does inflation targeting anchor long-run inflation ...swanson2/papers/termi.pdf · of the authors,...

DOES INFLATION TARGETING ANCHORLONG-RUN INFLATION EXPECTATIONS?EVIDENCE FROM THE U.S., UK, ANDSWEDEN

Refet S. GürkaynakBilkent University

Andrew LevinFederal Reserve Board

Eric SwansonFederal Reserve Bank of San Francisco

AbstractWe investigate the extent to which inflation expectations have been more firmly anchored inthe United Kingdom—a country with an explicit inflation target—than in the United States—acountry with no such target—using the difference between far-ahead forward rates on nominaland inflation-indexed bonds as a measure of compensation for expected inflation and inflationrisk at long horizons. We show that far-ahead forward inflation compensation in the U.S.exhibits substantial volatility, especially at low frequencies, and displays a highly significantdegree of sensitivity to economic news. Similar patterns are evident in the UK prior to 1997,when the Bank of England was not independent, but have been strikingly absent since the Bankof England gained independence in 1997. Our findings are further supported by comparisons ofdispersion in longer-run inflation expectations of professional forecasters and by evidence fromSweden, another inflation-targeting country with a relatively long history of inflation-indexedbonds. Our results support the view that an explicit and credible inflation target helps to anchorthe private sector’s views regarding the distribution of long-run inflation outcomes. (JEL: E31,E52, E58)

The editor in charge of this paper was Jordi Gali.Acknowledgments: In compiling the data for this project, we received invaluable help from JanAlsterlind, Andrew Clare, Lars Hörngren, Michael Joyce, Peter Lildholt, and Lena Stromberg. Thepaper has also benefited greatly from discussions, comments, and suggestions from Paul Beaudry,Alan Blinder, Mick Devereux, Ken Kuttner, Rick Mishkin, Scott Roger, Brian Sack, Klaus Schmidt-Hebbel, Eric Parrado, Lars Svensson, Jonathan Wright, and seminar participants at the SwedishRiksbank, the International Monetary Fund, Johns Hopkins University, UC Berkeley, the EuropeanCentral Bank, Università Bocconi/IGIER, Bilkent University, the European University Institute,the Federal Reserve Board, European Summer Symposium in Macroeconomics, and the NBERMonetary Economics Program Meeting. We appreciate the excellent research assistance of AndrewMarder, Claire Hausman, and Oliver Levine. The views expressed in this paper are solely thoseof the authors, and do not necessarily reflect the views of the Board of Governors of the FederalReserve System, the management of the Federal Reserve Bank of San Francisco, or any other personassociated with the Federal Reserve System.E-mail addresses: Gürkaynak: [email protected]; Levin: [email protected]; Swanson: [email protected]

Journal of the European Economic Association December 2010 8(6):1208–1242© 2010 by the European Economic Association

Gürkaynak et al. Does Inflation Targeting Anchor Expectations? 1209

1. Introduction

Long-term price stability is a central goal of monetary policy for virtually everymodern central bank.1 To facilitate the achievement of this objective, a numberof national and supranational central banks have adopted an “inflation target-ing” framework, in which a numerical objective for inflation is explicitly stated,vigorously pursued, and clearly communicated to the public in the form of peri-odic, detailed reports on the current and projected outlook for inflation and otheraspects of the macroeconomy (cf. Bernanke et al. 1999).2 The adoption of infla-tion targeting has been encouraged by a growing body of literature regarding theadvantages of this framework in the formulation and communication of monetarypolicy (e.g., Persson and Tabellini 1993; Walsh 1995).3

Nevertheless, empirical analysis using quarterly realizations of inflation orsurvey-based measures of inflation expectations has so far yielded at best weaksupport for the notion that inflation targeting (IT) significantly influences thebehavior of inflation. In particular, quarterly inflation rates and short-term inflationforecasts have not behaved very differently in IT and non-IT economies, becauseall of the major industrial nations experienced significant disinflation in the early-to-mid 1990s (Ball and Sheridan 2005; Gertler 2005).4 Moreover, analysis oflonger-term inflation expectations has been hampered by a scarcity of data dueto the relatively recent adoption of IT in most countries and the low (typicallysemiannual) frequency of surveys that measure long-run inflation expectations(Levin, Natalucci, and Piger 2003).

In this paper, we evaluate the influence of inflation targeting on long-terminflation expectations by comparing the behavior of daily bond yield data in theUnited States and the United Kingdom. Both countries have highly liquid marketsfor nominal and inflation-indexed government bonds across a wide range of matu-rities, which allows for the computation and comparison of forward nominal andreal interest rates in each country.5 Forward inflation compensation—defined

1. In other periods, of course, one can find many instances in which a central bank’s primaryobjective was to provide the government with cheap credit and seigniorage revenue.2. See also Leiderman and Svensson (1995), Bernanke and Mishkin (1997), and Kuttner (2005).3. See also Svensson (1997), McCallum (1996), Bernanke et al. (1999), and Svensson and Woodford(2003).4. See also Bernanke et al. (1999) and Johnson (2002).5. In ongoing research, we are working to extend the methods of this paper to other inflation-targeting countries. However, the data limitations for other countries are often severe or prohibitive:for example, New Zealand has only one inflation-indexed bond outstanding, which makes the com-putation of forward rates impossible. Canada had only one inflation-indexed bond until 1996 andonly two from 1996 to 2001, and even these bonds have extremely long durations (30 years) andlow liquidity, making implied forward rates difficult to estimate and noisy. High-frequency data onmarket forecasts of macroeconomic statistical releases in Australia, New Zealand, and Finland arenot available, to our knowledge. Finally, data in developing countries with inflation targets, such asSouth Africa and Chile, tends to be even more limited.

1210 Journal of the European Economic Association

as the difference between the forward rates on nominal and inflation-indexedbonds—measures the compensation that investors demand to cover the expectedrate of inflation and the risks associated with that inflation at a given horizon.6

In contrast to previous empirical studies of inflation targeting, the dailyfrequency of our bond yield data, together with the frequent release of impor-tant macroeconomic statistics and monetary policy announcements, enablesus to obtain relatively precise estimates of the impact of these news releaseson far-ahead forward inflation compensation, even for samples spanning onlyhalf a decade or so. If far-ahead forward inflation compensation is relativelystable and insensitive to incoming economic news, then that would suggestthat financial market participants have fairly stable views regarding the dis-tribution of long-term inflation outcomes, and hence that the monetary policyframework has been reasonably successful in anchoring long-term inflationexpectations.

Our analysis reveals substantial differences between the U.S. and UK withrespect to both the unconditional and conditional behavior of far-ahead forwardinflation compensation. For the United States, we find that far-ahead forward infla-tion compensation is quite volatile, especially at lower frequencies, and exhibitshighly significant responses to economic announcements. Furthermore, the mag-nitude of these responses does not diminish after a day or two, as one might haveexpected if economic news were merely inducing transitory fluctuations in mar-ket liquidity. Interestingly, we find very similar results for the UK prior to 1997,when the Bank of England was not independent, but not for the period since mid1997, when the Bank of England was independent. In particular, for the post-1997 sample of UK data, we find that far-ahead forward inflation compensationexhibits very little volatility, especially at low frequencies, and does not respondsignificantly to economic news. These results support the view that a transparentand credible inflation target helps to anchor the private sector’s perceptions of thedistribution of long-run inflation outcomes.

Importantly, our analysis of inflation compensation implicit in long-term bondyields does not rely on the expectations theory of the term structure; that is, thesepatterns need not be due solely to shifts in the conditional mean of the inflationrate at long horizons.7 In particular, if inflation expectations are not completelyanchored, then economic news might well shift the far-ahead forward inflationrisk premium, either because near-term economic developments affect investors’perceptions regarding the distribution of long-run inflation outcomes, or becausethe economic news has a significant impact on the price that investors attach

6. In contrast to yields, the use of forward rates avoids any direct influence from short-term devel-opments, thereby permitting a sharper focus on inflation expectations at a particular horizon. SeeSection 2.7. For empirical evidence regarding the failure of the expectations hypothesis, see Fama and Bliss(1987), Campbell and Shiller (1991), and Cochrane and Piazzesi (2005).


to those long-run inflation risks. In contrast, if inflation expectations are firmlyanchored, with a time invariant distribution around the specified target value, theneconomic news should have a much smaller impact on far-ahead forward inflationcompensation.

Our conclusions about the impact of inflation targeting are also bolsteredby some additional sources of evidence. First, we consider surveys of pro-fessional forecasters and document the extent to which dispersion in long-runinflation expectations is markedly higher for the United States than for the UnitedKingdom. Second, we analyze daily bond yield data from Sweden, anotherinflation-targeting country with a history of inflation-indexed government secu-rities spanning a range of maturities, albeit traded in markets that are notablyless liquid than those of the United Kingdom or the United States. Our resultsusing the Swedish data match those of the post-1997 UK data, namely, Swedisheconomic news does not have any significant influence on far-ahead forwardinflation compensation, consistent with the view that long-run inflation expecta-tions in Sweden are firmly anchored. Moreover, these results further underminethe notion that the U.S. findings might be due to fluctuations in market liquidityrather than to movements in expected inflation and perceived inflation risk at longhorizons.

The remainder of the paper proceeds as follows. Section 2 describes ourdaily data and how we compute forward nominal and real interest rates, inflationcompensation, and the surprise components of macroeconomic data releases andmonetary policy announcements. Section 3 compares the unconditional volatil-ities of far-ahead forward inflation compensation in the United States and theUnited Kingdom. Section 4 evaluates the extent to which U.S. far-ahead forwardinflation compensation is sensitive to economic news, and Section 5 performs thisanalysis using UK data over the pre-independence and post-independence sam-ple periods. Section 6 presents additional evidence from U.S. and UK surveys ofprofessional forecasters and from Swedish bond yield data. Section 7 concludes.An Appendix at the end of the paper provides further details regarding all of thedata series.

2. Methods and Data

If the steady-state inflation rate is constant over time and known by all agents—that is, if inflation expectations are well-anchored—then standard macroeconomicmodels predict that inflation should return to its steady state well within 10 yearsafter a shock (Gürkaynak, Sack, and Swanson 2005; Gürkaynak, Levin, and Swan-son 2006). To test whether this prediction is met in the data, we must look beyondthe effects of economic announcements on the first few years of the term structureand focus instead on the response of far-ahead forward interest rates and inflationcompensation to the announcement.


2.1. Forward Interest Rates and Forward Inflation Compensation

Forward rates are often a very useful means of interpreting the term structure ofinterest rates. For a bond with a maturity of m years, the yield r

(m)t represents

the rate of return that an investor requires to lend money today in return for asingle payment m years in the future (for the case of a zero-coupon bond). Bycomparison, the k-year-ahead one-year forward rate f

(k)t represents the rate of

return from period t +k to period t +k+1 that the same investor would require tocommit at time t to a one-year loan beginning at time t + k and maturing at timet + k + 1. The linkage between these concepts is simple: An m-year zero-couponsecurity can be viewed as a sequence of one-year forward agreements over thenext m years. The k-year-ahead one-year forward rate f

(k)t can thus be obtained

from the yield curve by the simple definition:8

1 + f(k)t =

(1 + r

(k+1)t

)k+1

(1 + r

(k)t

)k. (1)

For the U.S., we use data on nominal and real forward rates on U.S. Treasurysecurities produced by the Federal Reserve Board going back to 1998.9 For theUK, we use data on nominal and real forward rates on UK government securitiesproduced by the Bank of England going back to 1985.10 For Sweden, we com-puted nominal and real forward rates from data on nominal and inflation-indexedSwedish government yields obtained from the Swedish Riksbank going back

8. If we observed zero-coupon yields directly, computing forward rates would be as simple as this.In practice, however, most government bonds make regular coupon payments and thus the size andtiming of the coupons must be accounted for to translate observed yields into the implied zero-couponyield curve. In Gürkaynak, Levin, and Swanson (2006), we also reported results using U.S. TreasurySTRIPS data, which are actually traded zero-coupon securities, and showed that our results are notsensitive to using actually traded as opposed to implied zero-coupon yields. Note also that our yieldcurve data for the U.S., UK, and Sweden are all quoted on a continuously compounded basis, whichimplies that our forward rate data is given by f

(k)t = (k + 1)r

(k+1)t − kr

(k)t rather than equation (1),

which is for annually compounded yields.9. The Federal Reserve Board computes daily implied zero-coupon yields from off-the-run U.S.Treasury yields using the extension of the Nelson–Siegel (1987) method proposed by Svensson(1994); see Gürkaynak, Sack, and Wright (2005) for details. U.S. inflation-indexed bonds (TIPS)were issued for the first time in January 1997 and only annually in the first few years after that date,so our far-ahead forward real rate data for the U.S. begins in January 1998. Although the FederalReserve Board provides a full estimated real yield curve only back to January 1999 (Gürkaynak,Sack, and Wright, 2010), we extend the 9–10 year forward rate series back to January 1998 by takingthe 9- and 10-year TIPS rates and computing the implied forward rate between the two using theShiller–Campbell–Schoenholtz (1983) approximation.10. The Bank of England computes implied zero-coupon nominal and real yields from observedUK government yields using a spline-based procedure (details and data are available from the Bankof England’s Web site), with daily data available back to 1985.


to 1996.11 Although the yield curves of different countries are estimated usingsomewhat different methodologies, this has no effect on our results as all of theseare very well-fitting yield curves that reveal the underlying discount functionsappropriately. Having obtained forward nominal rates and forward real rates foreach country, we define forward inflation compensation to be the forward nominalrate less the forward real rate at each horizon. Note that this measure captures thecompensation that investors demand both for expected inflation and for the risksor uncertainty associated with that inflation at that horizon.

Given our interest in measuring long-term expectations, we focus our analysison the longest maturity for which we have high-quality data for both real andnominal bond yields. The exceptional liquidity, depth, and breadth of the marketsfor government securities near the 10-year horizon thus suggests focusing onthe one-year forward rate from 9 to 10 years ahead (i.e., the one-year forwardrate ending in 10 years). This horizon is sufficiently far out for standard macro-economic models to return essentially to steady state, so that any movements inforward inflation compensation at these horizons are very difficult to attribute totransitory responses of the economy to a shock.

2.2. Regression Specification

We compare the unconditional volatility of far-ahead forward inflation compensa-tion in the U.S. and UK and also the sensitivity of forward inflation compensationin each country to major macroeconomic announcements. To study the sensi-tivity of inflation compensation, we run a series of high-frequency event-studyregressions of the form

�yt = α + βXt + εt , (2)

where t indexes days, �yt is the change in forward inflation compensation overthe day, Xt is the surprise component of the macroeconomic data releases andmonetary policy announcements that took place that day, and εt is a residualrepresenting the influence of other factors on yt that day. As a benchmark forcomparison, we also run regressions of the form (2) with the change in the one-yearspot nominal interest rate on the left-hand side.

We now describe in detail the data that underlie this analysis.

11. For Sweden, we backed out the implied zero-coupon yield curves and forward rates usingthe Svensson (1994) methodology (which was designed for Swedish data and is the same methodemployed by the Federal Reserve Board for U.S. data) and checked that these did in fact fit theSwedish bond data very well. The first inflation-indexed Swedish government bond was issued inMarch 1994, but additional indexed bonds were not issued until May 1996, when a range of fournew maturities were issued, so our forward real rate data for Sweden begin in May 1996.


2.3. Macroeconomic Data Releases

Financial markets are forward-looking, so the expected component of macro-economic data releases should have essentially no effect on interest rates.12 Tomeasure the effects of macroeconomic data releases on interest rates, then, wemust first compute the unexpected, or surprise, component of each release. Usingthe surprise component of the releases also removes any issues of endogeneityarising from interest rates feeding back to the macroeconomy, because any sucheffects, to the extent that they are predictable, will be incorporated into marketexpectations for the release.

To measure the surprise component of each data release in our sample, wecompute the difference between the actual release and the median forecast of thatrelease made by professional forecasters just a few days prior to the event. Forthe U.S. and UK, we obtained data on professional forecasts of the next week’sstatistical releases for around fifty macroeconomic time series for each countrycollected and published every Friday by Money Market Services (MMS).13 How-ever, not all of these statistics have a significant impact on interest rates, even atthe short end of the yield curve. Thus, to conserve space and reduce the number ofexogenous variables in our regressions, we restrict attention to only those macro-economic variables that have the largest and most statistically significant effectson the spot one-year Treasury bill rate in those countries.14 Note that this selec-tion procedure does not bias our estimates of the sensitivity of far-ahead forwardinterest rates to economic news because those interest rates are insensitive to allmacroeconomic and monetary policy announcements under the null hypothesis.Our results herein are very similar when we include all available variables on theright-hand side of the regression.

In contrast to the U.S. and UK, data on professional forecasts is more limitedfor Sweden. Thus, for that country we use all of the available professional forecastdata collected by Bloomberg Financial Services every week.15

12. Kuttner (2001) tests and confirms this hypothesis for the case of monetary policy announce-ments.13. The quality of the MMS data as measures of expectations has been verified by previousauthors—see, for example, Balduzzi, Elton, and Green (2001) and Andersen et al. (2003).14. In particular, for the U.S. we begin with all releases and choose those that have a statisticallysignificant effect on the 1-year nominal rate. Core CPI is added to this list because this variableis highly significant in longer samples and belongs in a study of inflation expectations a priori.New home sales, another important release, did not make it into the final list. Its inclusion wouldhave made our results even stronger. For the UK and Sweden the variable lists are determined byavailability and our desire to span releases about prices, real activity and monetary policy.15. Bloomberg offers forecasts of a greater number of Swedish statistics than Money MarketServices and were more readily available to us. For the U.S. and UK, the Bloomberg forecast datado not go back as far as the MMS data (about 1996 for Bloomberg vs. 1985 for MMS-U.S. and 1993for MMS-UK), but the two data sources agree very closely when they overlap.


Additional details regarding the macroeconomic data releases and profes-sional forecast data for all of these countries are included in the Data Appendixat the end of this paper.

2.4. Monetary Policy Announcements

As with macroeconomic data releases, we must compute the surprise compo-nent of monetary policy announcements in each country in order to measure theeffects of these announcements on interest rates. Rather than use the medianof professional forecasts to measure expectations, however, we use the one-daychange in a short-term interest rate, such as a 3-month government bill rate,around each monetary policy announcement to measure the surprise componentof the announcement. The advantage of using market-based measures of mone-tary policy surprises is that they are of higher quality and are available essentiallycontinuously—see, for example, Krueger and Kuttner (1996), Rudebusch (1998),and Gürkaynak, Sack, and Swanson (2007).

For the U.S., we measure monetary policy surprises using the change inthe current-month federal funds future contract on the dates of Federal Reservemonetary policy announcements, as in Kuttner (2001). For the UK and Sweden,we do not have futures data for the policy rates of the corresponding centralbanks, so we measure monetary policy surprises using the change in the spot3-month UK government bill rate on the days of Bank of England monetarypolicy announcements and the change in the 3-month Swedish government billrate on the days of Riksbank monetary policy announcements. The change in the3-month rate on these days reflects changes in financial market expectations aboutthe current and future course of monetary policy over the subsequent 3 months—although this is not the same as the shorter horizon one would obtain from a verynear-term futures contract, it is nonetheless an excellent measure of the change inthe near-term monetary policy environment.16 Additional details regarding theseannouncements and financial market measures of the surprise component of theseannouncements are provided in the Data Appendix.

3. The Unconditional Volatility of Forward Inflation Compensation

3.1. The United Kingdom

Figure 1 illustrates the use of far-ahead forward inflation compensation to assessthe anchoring of long-term inflation expectations in the UK. In Figure 1, we plot

16. We have verified that using the 1-day change in the 3-month U.S. Treasury bill rate to measurethe surprise component of U.S. monetary policy announcements does not alter our findings. We stickwith federal funds futures as the preferred measure for the U.S. because that is the most commonin the literature and because Gürkaynak, Sack, and Swanson (2007) showed that, among the manypossible financial market instruments that potentially measure U.S. monetary policy expectations,federal funds futures are the most accurate in a forecasting sense.


Figure 1. The evolution of far-ahead forward inflation compensation in the United Kingdom, 1990–2005. The figure depicts the daily series of the nine-year-ahead one-year forward rate of inflationcompensation for the United Kingdom over the period 2 January 1990 through 28 December 2005.

the daily time series of 1-year forward UK inflation compensation 9 years ahead(that is, the forward nominal interest rate from 9 to 10 years ahead less the forwardreal rate from 9 to 10 years ahead). There are three particularly noteworthy eventsin the figure. First, in September 1992, this measure of inflation compensationsoared about 250 basis points over the course of a week, as the British governmentdropped out of the European Monetary System and adopted a floating exchangerate regime. Interestingly, the following month’s announcement of the adoption ofa 2.5% inflation target did not generate any noticeable decline in forward inflationcompensation, presumably reflecting the extent to which the announced inflationtarget was not viewed by financial markets as being particularly credible (a themeto which we will return subsequently).

Second, in early May 1997, Chancellor of the Exchequer Gordon Brown madea surprise announcement that the Bank of England would be granted operationalindependence from the Exchequer and Parliament.17 Far-ahead forward infla-tion compensation plummeted by an amazing 75 basis points that day, and thendeclined further during the remainder of the year as the new policy regime becameinstitutionalized and the Bank of England released its first Inflation Report. Fromearly 1998 through the end of our sample in 2005, far-ahead forward inflation

17. The 6 May 1997 announcement came as a surprise to financial markets, particularly the scopeof the announcement: According to the BBC, the “surprise announcement . . . is being described asthe most radical shake-up in the Bank’s 300-year history” (British Broadcasting Corporation 1997).Central bank independence was officially passed into law on 23 April 1998 with an effective date of1 June 1998.


compensation remained in the range of about 2% to 3%, apparently reflecting theconfidence of financial markets the Bank of England would indeed keep inflationclose to the prescribed target.

Finally, in early December 2003, the Chancellor of the Exchequer announcedthat the inflation target would be specified in terms of the consumer price index(CPI) instead of the retail price index excluding mortgage interest (RPIX), and thatthe numerical target for CPI inflation would be set at 2%, a half percentage pointlower than the previous target for RPIX inflation.18 However, as Bank of EnglandGovernor Mervyn King (2004) noted in a subsequent speech, this methodologicalchange effectively raised the inflation target by a bit less than half a percentagepoint, because RPIX inflation had typically exceeded CPI inflation by nearly a fullpercentage point over the previous decade. As evident from Figure 1, far-aheadforward inflation compensation rose by about 40 basis points in the wake of theadjustment to the inflation targeting regime and remained at this somewhat higherplateau throughout 2004 and 2005.

3.2. The United States

Figure 2 compares the behavior of far-ahead forward inflation compensation forthe United States with that of the United Kingdom.19 The differences betweenthe two series are striking, especially since the beginning of the current decade.First, the U.S. series has an average value close to 3%, more than 30 basis pointshigher than the average for the UK series. Second, the U.S. series has exhibitedmuch greater volatility, with an unconditional standard deviation of about 0.4%,roughly twice as high as the unconditional volatility of the UK series. Finally,the fluctuations in U.S. far-ahead forward inflation compensation are much morepersistent than in the UK data.

Figure 3 compares the spectrum of far-forward inflation compensation forthe United States with that of the United Kingdom.20 Spectral analysis provides aconvenient way of visualizing the persistence of each series, because the uncon-ditional variance is given by the integral of the spectrum, and the height of thespectrum at each frequency shows the extent to which stochastic fluctuationsat that frequency contribute to the overall variance. As is evident in the figure,

18. The CPI, or Harmonized Index of Consumer Prices (HICP), was regarded as a better measureof inflation than the core Retail Price Index (RPI), which included some interest financing costs,among other things. Moreover, the HICP was the standard measure of inflation being adopted bymembers of the European Monetary Union. Inflation indexed government securities in the UK areindexed to the RPI.19. Starting in January 2004, the UK series incorporates a constant 40-basis-point adjustment thatreflects the switch from RPIX to CPI in the definition of the inflation target.20. The spectrum is computed using autoregressive spectral density estimation (cf. Priestley1981).


Figure 2. Comparing the evolution of far-ahead forward inflation compensation in the UnitedKingdom vs. the United States, 1998–2005. This figure depicts the daily series of 9-year-ahead1-year forward rates of inflation compensation for the United States (dashed line) and the UnitedKingdom (solid line) over the period 2 January 1998 through 29 December 2005. Starting in January2004, the UK series incorporates a constant 40-basis-point adjustment that reflects the switch fromRPIX to HICP in the definition of the inflation target.

Figure 3. The spectral decomposition of far-ahead forward inflation compensation in the UnitedKingdom vs. the United States, 1998–2005. This figure depicts the spectral density function of the9-year-ahead 1-year forward rate of inflation compensation for the United States (dashed line) and forthe United Kingdom (solid line), using monthly average data for 1998:01 through 2005:12. Startingin January 2004, the UK series incorporates a constant 40 basis point adjustment that reflects theswitch from RPIX to HICP in the definition of the inflation target. For each series, the spectral densityis shown over frequencies from 0 to π/2, corresponding to cycles lasting 2 months or longer.


U.S. and UK forward inflation compensation exhibit roughly similar magnitudesof high-frequency, transitory variation of the sort that might be associated withfluctuations in market liquidity or other technical factors.

In contrast, the persistent component of the volatility of the U.S. data isdramatically higher than that of the UK data, fully accounting for the differ-ence between the unconditional variances of the two series. These unconditionalmoments of the data foreshadow the key results that we obtain below, namely,that the greater volatility of U.S. far-ahead forward inflation compensation seemsto arise from systematic and persistent movements in investors’ inflation expec-tations (and perceived inflation risk at long horizons) and not from transitorytechnical factors or market noise.

4. The Sensitivity of U.S. Inflation Compensation to Economic News

Although far-ahead forward inflation compensation in the U.S. has been morevolatile than in the UK, especially with respect to the persistent component ofthe two series, it is natural to ask whether and to what extent this difference hasbeen systematic. That is, does U.S. inflation compensation appear to fluctuaterandomly, perhaps because of liquidity or other technical factors, or does U.S.inflation compensation appear to respond systematically to major macroeconomicnews? A systematic difference in responsiveness would suggest that the highervolatility of U.S. inflation compensation is not due to liquidity or other technicalbond market behavior, but is instead related to varying bond market expectationsor concerns regarding the distribution of long-run inflation outcomes.

Table 1 reports the results for regression (2) applied to far-ahead forwardinflation compensation and also to the spot one-year nominal interest rates asa benchmark for comparison. Each of the two columns reports the results froma regression of daily changes in the short-term interest rate (or far-ahead for-ward inflation compensation) on the surprise component of the major economicannouncements listed at the left. We restrict attention in the regressions to onlythose days on which some macroeconomic statistic was released or a monetarypolicy announcement was made, but our results are not sensitive to this restriction.Note that, although there are nearly 800 daily observations in each of these regres-sions on which some major economic announcement was made, most of thoseobservations for any individual regressor are zero because any given macroeco-nomic statistic is only released once per month (or once per quarter in the case ofGDP, once per week in the case of Initial Claims).

To aid in interpreting our coefficient estimates, each macroeconomic surpriseis normalized by its standard deviation, so that the coefficients in the table reportthe interest rate response in basis points per standard deviation surprise in thecorresponding macroeconomic statistic—the one exception to this rule is for


Table 1. U.S. forward rate responses to economic news (1998–2005).

1-year Forward Inflation1-year Compensation

Nominal Rate ending in 10 yrs

Core Consumer Price Index 1.23 1.51∗(1.74) (2.29)

Real GDP (advance) 2.18∗ 1.77∗(2.50) (2.06)

Initial Jobless Claims −1.10∗∗ −0.49∗(−3.72) (−2.03)

NAPM/ISM Manufacturing 2.29∗∗ 1.48∗(2.66) (2.56)

New Home Sales 0.62 1.44∗∗(1.59) (3.50)

Nonfarm Payrolls4.54∗∗ 0.54(7.47) (0.84)

Monetary Policy 0.24∗ −0.12(2.07) (−1.42)

# Observations 787 787R2 .14 .04Joint test p-value .0000∗∗ .0000∗∗

Notes: Sample: Jan 1998–Dec 2005 at daily frequency on the dates of macroeconomic and monetary policyannouncements. Heteroskedasticity-consistent t-statistics reported in parentheses. ∗Significant at 5%; ∗∗Significant at1%. Regressions also include a constant, a Y2K dummy that takes on the value 1 on the first business day of 2000, and ayear-end dummy that takes on the value 1 on the first business day of any year (coefficients not reported). Macroeconomicdata release surprises are normalized by their standard deviations, so that coefficients represent a basis point per standarddeviation response. Monetary policy surprises are in basis points, so that those coefficients represent a basis point perbasis point response. Inflation compensation is the difference between nominal and real rates. Joint test p-value is for thehypothesis that all seven coefficients (other than the constant and dummy variables) are zero. See text for details.

monetary policy surprises, which we leave in basis points, so that those coefficientsrepresent a basis-point per basis-point response.21

4.1. The Impact of News on U.S. Short-Term Treasury Rates

The first column of Table 1 reports the responsiveness of the spot one-year U.S.Treasury rate to our macroeconomic and monetary policy announcements as abenchmark for the far-ahead forward inflation compensation results. Not surpris-ingly, we find that this short-term interest rate responds to these announcementswith an overwhelming degree of statistical significance—the p-value for the jointhypothesis that all of the coefficients are equal to zero, excepting the constant anddummy variables, is below 10−15. Moreover, the responses we estimate are all

21. Each regression also includes a constant, a “Y2K” dummy that takes on the value 1 on the firstbusiness day of 2000, and a year-end dummy that takes on the value 1 on the first business day ofany year. These coefficients are not reported to save space and because we generally found them tobe unimportant in our regressions.


consistent with what one would expect from a Taylor-type reaction function formonetary policy: Upward surprises in inflation, output, or employment lead toincreases in short-term interest rates, and upward surprises in initial jobless claims(a countercyclical economic indicator) causes short-term interest rates to fall.

Although the magnitudes of the response coefficients in Table 1 might seemsmall at first glance—a two-standard-deviation surprise leads on average to a3-basis-point change in the 1-year rate—they are in fact not surprising given therelatively high noise-to-signal ratio of the monthly data releases for the true under-lying level of economic activity and rate of inflation. Similarly, the regression R2

is only about 14%, implying that even on those 787 days when we know the majoreconomic news that day, the regression explains only one-seventh of the variationin short-term rates due to the complexities of the announcement—our surprisedata is only for the headline component of the release, whereas the full release isoften much richer22—and other factors influencing Treasury yields. Nevertheless,the extraordinary statistical significance of many of the individual coefficients andthe regression as a whole imply that the economic releases in Table 1 do containinformation that is extremely relevant for the behavior of short-term interest rates.

A final point worth noting in the short-rate regression in Table 1 is thatmonetary policy surprises lead to about a 1-for-4 response of the one-year yieldto the federal funds rate. This is consistent with the view that a surprise change inthe funds rate is generally not a complete surprise to markets, but rather a bringingforward or pushing back of policy changes that were largely expected to occurwithin the next year, anyway.23

4.2. The Impact of News on U.S. Far-Forward Inflation Compensation

The second column of Table 1 addresses the central question of the paper: Doesfar-ahead forward inflation compensation in the U.S. respond systematically toeconomic news? If 10 years is a sufficient length of time for U.S. inflation toreturn essentially to steady state following an economic shock, and if long-terminflation expectations are firmly anchored in the U.S., then we would expectto see no systematic response of far-ahead forward inflation compensation toeconomic news. As is apparent in the second column of Table 1, this is not thecase: Ten-year-ahead forward inflation compensation responds significantly—and often very highly so—to five of the seven macroeconomic announcements in

22. For example, the CPI release includes not just the top-line inflation numbers, but also a detailedbreakdown of inflation by product category. Markets may respond differently to a given headlinenumber depending on the underlying detail of the release and whether that detail suggests the newsin the release is transitory or more permanent. Our professional forecast data covers only the top-linenumber. The situation is very similar for GDP and indeed all of the other macroeconomic statisticswe consider.23. Gürkaynak (2005) discusses this point in more detail.


the table, and the joint hypothesis that all of the coefficients in the regression arezero is rejected with a p-value below one in ten million.24 Moreover, the signsand magnitudes of these coefficients are not random, but rather closely resemblethose from the first column—that is, positive surprises to output or inflation causefar-ahead forward inflation compensation to rise in line with short-term rates. Thisis precisely the result we would expect if financial markets expected some degreeof (or some chance of) pass-through from short-term inflation to the long-terminflation outlook.

The one exception to this empirical pattern is monetary policy announce-ments, for which the estimated effect on far-ahead forward inflation compensationis opposite the effect on the spot 1-year rate—that is, a surprise monetary policytightening causes far-ahead forward inflation compensation to fall. Although thiseffect is not statistically significant in Table 1, Gürkaynak, Sack, and Swanson(2005) show that this same effect is statistically significant for far-ahead forwardnominal interest rates over sample periods that extend back to 1990 or 1994.Again, this pattern is exactly what we would expect if financial markets expectedsome degree of (or some chance of) pass-through from short-term inflation to thelong-term inflation outlook.25

As was the case in the first column, the regression R2 and magnitudes of thecoefficients in the second column of Table 1 might seem small at first glance.However, the R2 in the second column is about one-third the size of that for theone-year Treasury rate in the first column, suggesting a large degree of explanatorypower for an interest rate which, under the null, should be a constant or statisticalwhite noise. Moreover, the sensitivity of inflation compensation to economic newsin the second column is almost as large as, and has the same sign as, the sensitivityof the short-term interest rate to these announcements. Because we know thatshort-term interest rates respond importantly to news about output and inflation,the sensitivity of far-ahead forward inflation compensation in the second columnshould also be regarded as being economically as well as statistically significant.Finally, although the effect of any single monthly announcement is only a fewbasis points, the effects of these announcements cumulate across releases andover time. Thus, the few basis points per announcement that we estimate oftenadd up, over the course of just a few months, to large and significant changes inlong-term interest rates and inflation compensation.

24. In the working version of this paper (Gürkaynak, Levin, and Swanson 2006), we also reportedresults for far-ahead forward nominal and real rates separately, but we do not report those here in theinterests of space. By and large, far-ahead forward real rates appear unresponsive to data surprisesin our sample. The signs of coefficients are mixed and there are very few significant coefficients.Beechey and Wright (2008) carry out a detailed study of the response of real rates.25. Gürkaynak, Sack, and Swanson (2005) present a simple New Keynesian model which hasthe feature that long-term inflation expectations, rather than being perfectly anchored, exhibit somedegree of loading on the recent history of inflation, and show that this model is consistent with allof the empirical findings in Table 1.


Figure 4. Daily changes in U.S. far-ahead forward inflation compensation. The left panel denotesquarterly observations of the percentage point surprise in the advance release of U.S. GDP (nor-malized by the standard deviation of these surprises) and the corresponding 1-day change in the9-year-ahead 1-year forward rate of inflation compensation (in basis points); the right panel pro-vides similar information for surprises in the monthly release of the U.S. consumer price indexexcluding food and energy. In each panel, the solid line indicates the least-squares regression line.

4.3. Sensitivity Analysis

Before drawing definitive conclusions from these regression results, some sensi-tivity analysis is certainly merited. We have performed a battery of outlier tests,which confirm that neither the magnitude of the estimated coefficients nor theirstatistical significance is sensitive to the exclusion of any single observation orpair of observations. The robustness of the results is also evident from the scatterplots in Figure 4. Each point in the left panel denotes the surprise in the advancerelease of GDP (horizontal axis) plotted against the corresponding one-day changein far-ahead forward inflation compensation (vertical axis), and the right panelprovides similar information for core CPI surprises; in both cases, the systematicresponse of far-ahead forward inflation compensation is readily apparent.

We also consider whether the impact of economic news on far-ahead forwardinflation compensation might be purely transitory, as one might expect if thesesystematic effects mainly reflected fluctuations in market liquidity or varioustechnical factors. The first column of Table 2 repeats the regression results fromthe previous table (that is, the same-day response of inflation compensation tomacroeconomic and monetary policy surprises), and the second and third columnsreport parallel results where the dependent variable is either the 2-day change orthe 3-day change in far-ahead forward inflation compensation.

With the exception of surprises in the NAPM survey, these results generallyindicate that these surprises have persistent effects on far-ahead forward inflation


Table 2. Multiday responses of U.S. long-run forward inflation compensation to economicnews (1998–2005).

Same-Day Response After Response AfterResponse One Day Two Days

Core Consumer Price Index 1.51∗ 2.06∗ 1.44(2.29) (2.21) (1.51)

Real GDP (advance)1.77∗ 2.29∗ 2.71(2.06) (2.16) (1.88)

Initial Jobless Claims −0.49∗ −0.54 −0.75∗(−2.03) (−1.36) (−1.97)

NAPM/ISM Manufacturing 1.48∗ 0.57 0.40(2.56) (0.71) (0.56)

New Home Sales 1.44∗∗ 1.38∗ 2.18∗∗(3.50) (2.26) (3.16)

Nonfarm Payrolls0.54 1.44 1.38

(0.83) (1.42) (1.41)

Monetary Policy −0.12 −0.08 −0.06(−1.42) (−0.75) (−0.41)

# Observations 787 971 967R2 .04 .02 .03Joint test p-value .0000∗∗ .02∗ .01∗∗

Notes: Sample: Jan 1998–Dec 2005 at daily frequency on the dates of macroeconomic and monetary policyannouncements. Heteroskedasticity-consistent t-statistics reported in parentheses. ∗Significant at 5%; ∗∗Significant at1%; Regressions also include a constant, a Y2K dummy that takes on the value 1 on the first business day of 2000, and ayear-end dummy that takes on the value 1 on the first business day of any year (coefficients not reported). Macroeconomicdata release surprises are normalized by their standard deviations, so that coefficients represent a basis point per standarddeviation response. Monetary policy surprises are in basis points, so that those coefficients represent a basis point perbasis point response. Inflation compensation is the difference between nominal and real rates. Joint test p-value is for thehypothesis that all seven coefficients (other than the constant and dummy variables) are zero. See text for details.

compensation, consistent with the view that this news has a systematic influenceon long-run inflation expectations and inflation risks. For the core CPI surprises,the magnitude of the regression coefficient is similar in all three columns ofTable 2, confirming that the initial impact on far-ahead forward inflation com-pensation is not reversed over the next several days. For several other explanatoryvariables (namely, surprises in advance GDP, initial jobless claims, new homesales, and nonfarm payrolls), the estimated impact on far-ahead forward inflationcompensation even grows over the course of 2 or 3 days; these patterns could indi-cate that the same-day effects of these surprises tend to be dampened by liquidityeffects, or alternatively could reflect the extent to which some economic newsreceives further attention from market commentators and then induces furtherreverberations on inflation compensation.

5. The Sensitivity of UK Inflation Compensation to Economic News

Given the previous evidence that U.S. far-ahead forward inflation compensationhas been more sensitive to economic news than one would expect if inflationexpectations were perfectly anchored, we now investigate the extent to which


similar relationships are apparent in UK data. To shed further light on the role ofthe monetary policy regime, we first consider the sample period prior to mid 1997and then turn to the period over which the Bank of England has had operationalindependence from the Government.

5.1. The Period Prior to Central Bank Independence

Although the United Kingdom officially adopted an inflation target in October1992, the interest rate policy of the Bank of England continued to be set by theChancellor of the Exchequer, who was a member of Parliament and of the PrimeMinister’s Cabinet. The lack of operational independence of the Bank of Englandarguably led to a diminished level of credibility, and survey data as well as bondyields indicate that long-term inflation expectations remained substantially higherthan the official inflation target of 2.5%.26

Table 3 reports regression results for UK data over the sample period fromFebruary 1993 to April 1997.27 The format of the table is the same as Table 1, andindeed, the results here are quite similar to those for the United States. In particular,for this period, UK far-ahead forward inflation compensation appears to be verysensitive to economic news: Four of the seven coefficients exhibit a high degreeof statistical significance, and the p-value indicates an overwhelming rejection ofthe joint hypothesis that these announcements have no systematic impact on thedependent variable. As in the U.S., these coefficients are very similar in magnitudeto those for short-term rates, implying that the sensitivity is economically as wellas statistically significant. Indeed, UK far-ahead forward rates during this periodare even more sensitive to economic news, as measured by the regression R2,both in absolute terms and relative to short-term interest rates.

Interestingly, UK far-ahead forward inflation compensation respondedinversely to monetary policy surprises over this period, to an extent that is evengreater than was the case for the United States; that is, the UK estimates indicatethat an unexpected easing or tightening of the stance of monetary policy wouldtend to have an even greater impact on far-ahead forward inflation compensa-tion. This brings us to our final observation, which is that all of the significantcoefficients in Table 3, including those on monetary policy announcements, areconsistent with the view that during the period preceding the operational inde-pendence of the Bank of England, UK financial markets expected that near-termeconomic and monetary policy surprises would have substantial effects on thedistribution of inflation outcomes at very long horizons.

26. See Section 6.1 herein and Figure 1.27. Recall that the MMS data for the UK begin in February 1993. Moreover, there is a change inUK exchange rate and monetary policy regime in September 1992 which also suggests beginningthe sample in January 1993.


Table 3. UK forward rate responses to economic news, pre–Bank of England independence(1993–April 1997).

1-year Forward Inflation1-year Nominal Rate Compensation ending in 10 yrs

Average Earnings 3.23∗∗ 0.15(3.33) (0.20)

Real GDP (preliminary) 1.75 1.80∗(1.68) (2.02)

Manufacturing Production 0.76 0.33(0.88) (0.29)

Producer Price Index 2.13∗∗ 2.22∗(3.12) (2.61)

Core Retail Price Index 2.39∗∗ 2.60∗∗(3.19) (3.08)

Retail Sales2.17∗∗ −0.19(2.98) (−0.24)

Monetary Policy 0.67∗∗ −0.60∗∗(5.73) (−5.99)

# Observations 237 237R2 .35 .21Joint test p-value .0000∗∗ .0000∗∗

Notes: Sample is Feb 1993–Apr 1997 at daily frequency on the dates of macroeconomic and monetary policyannouncements. Heteroskedasticity-consistent t-statistics reported in parentheses. ∗Significant at 5%; ∗∗Significant at1%. Regressions also include a constant and a year-end dummy that takes on the value 1 on the first business day ofany year (coefficients not reported). Macroeconomic data release surprises are normalized by their full-sample standarddeviations, so that coefficients represent a basis point per standard deviation response. Monetary policy surprises arein basis points, so that those coefficients represent a basis point per basis point response. Inflation compensation is thedifference between nominal and real rates. Joint test p-value is for the hypothesis that all seven coefficients (other thanthe constant and dummy variables) are zero. See text for details.

5.2. The Period Following Central Bank Independence

On 6 May 1997, Chancellor of the Exchequer Gordon Brown announced that theBank of England would be granted independence to set its own interest rate policy,and this motion was passed into law by Parliament on 23 April 1998, with aneffective date of 1 June 1998. Thus, using a sample from July 1998 to December2005, we now consider whether UK far-ahead forward inflation compensationcontinued to remain sensitive to economic news even after the Bank of Englandgained operational independence.

Indeed, as shown in Table 4, the effects of news are strikingly different for thepost-independence sample: Although short-term interest rates in the first columncontinue to respond to economic news in very much the same way as in theearlier period, far-ahead forward inflation compensation now shows essentiallyno sensitivity to economic news. The series of retail sales surprises is the onlyexplanatory variable with a statistically significant impact, and this variable has anegative coefficient that is inconsistent with the notion that near-term inflationaryeffects might pass through to the longer-term inflation outlook. Moreover, the jointhypothesis of zero coefficients on all explanatory variables cannot be rejected at


Table 4. UK forward rate responses to economic news, post-Bank of England independence(July 1998–2005).


Average Earnings 1.81∗∗ −0.26(4.12) (−0.94)

Real GDP (preliminary) 2.04∗∗ −0.49(4.02) (−0.54)

Manufacturing Production 1.26∗∗ −0.04(3.09) (−0.09)

Producer Price Index 0.21 −0.22(0.55) (−0.63)

Core Retail Price Index 2.60∗∗ −0.76(4.83) (−1.95)

1.58∗∗ −1.18∗∗Retail Sales (3.92) (−2.75)

Monetary Policy 0.72∗∗ −0.13(5.96) (−1.01)

# Observations 480 480R2 .24 .03Joint test p-value .0000∗∗ .051

Notes: Sample: July 1998–Dec 2005 at daily frequency on the dates of macroeconomic and monetary policyannouncements. Heteroskedasticity-consistent t-statistics reported in parentheses. ∗Significant at 5%; ∗∗Significant at1%. Regressions also include a constant, a Y2K dummy that takes on the value 1 on the first business day of 2000, and ayear-end dummy that takes on the value 1 on the first business day of any year (coefficients not reported). Macroeconomicdata release surprises are normalized by their full-sample standard deviations, so that coefficients represent a basis pointper standard deviation response. Monetary policy surprises are in basis points, so that those coefficients represent a basispoint per basis point response. Inflation compensation is the difference between nominal and real rates. Joint test p-valueis for the hypothesis that all seven coefficients (other than the constant and dummy variables) are zero. See text for details.

standard significance levels, and the regression R2 is small, both in absolute termsand relative to the explainable variation in short-term interest rates. Thus, a knownand credible inflation target seems to have anchored the perceived distribution oflong-run inflation outcomes in the UK.

5.3. Sensitivity Analysis

As for the U.S. analysis, we have performed a battery of outlier tests to confirmthat the UK results are not sensitive to the exclusion of any single observationor pair of observations. The robustness of these results is also evident from thescatter plots in Figure 5, where the left column of panels provides informationabout the pre-independence sample (February 1993 to April 1997), and the rightcolumn of panels gives parallel information for the post-independence sample(July 1998 to December 2005). Each scatter plot depicts the 1-day response offar-ahead forward inflation compensation to surprises in the advance release ofreal GDP (A), surprises in the core consumer price index (B), and surprises inmonetary policy announcements (C).


Figure 5. Daily changes in UK far-ahead forward inflation compensation. Each panel depicts the1-day change in the UK 9-year-ahead 1-year forward rate of inflation compensation (in basis points)in response to a percentage point surprise in the advance release of UK GDP (A), the consumer priceindex (B), and monetary policy announcements (C), where each surprise has been normalized bythe standard deviation of that series of surprises. For each panel, the left panel depicts data for thepre-independence sample period (February 1994 to May 1997), and the right panel depicts data forthe post-independence sample period (July 1998 to December 2005). In each panel, the solid lineindicates the least-squares regression line.


The contrast between the two sample periods is readily apparent: Thesystematic influence of economic and monetary policy surprises during thepre-independence period is completely absent from the post-independence data.Moreover, one specific outlier—the huge positive monetary policy surprise in thelower-right panel—is particularly noteworthy. On that day, 12 September 1994,Chancellor of the Exchequer Kenneth Clarke

became the first chancellor in living memory to take the unpopular step ofraising interest rates not in response to soaring prices or a sterling crisis, butas a prudent move against future inflation… . Financial markets have hithertobeen sceptical of the government’s ability to meet its inflation target… . Thechancellor’s display of mettle strengthened his government’s credibility and,as a result, caused long-term interest rates to fall. (The Economist, 1994)

Several aspects of the narrative in The Economist are consistent with the viewthat UK long-term inflation expectations were not very well anchored at that time.First, this datapoint represents a genuine surprise, not just an anomaly in the data,and hence bolsters the strong statistical significance of the coefficient on monetarypolicy announcements in that regression. Second, despite the existence of anofficial inflation target prior to Bank of England independence, the Economiststory emphasizes that financial markets were skeptical about the Chancellor’scommitment to the inflation target. Finally, the article directly attributes that day’smovement in UK long-term interest rates as a response to the surprise in monetarypolicy, and, in particular, the extent to which this surprise influenced investors’perceptions of the long-term inflation outlook—precisely the explanation thatseems consistent with all of our findings.28

6. Additional Evidence

All of these results are consistent with the hypothesis that financial market percep-tions of the distribution of long-run inflation outcomes in the U.S. and pre-1997UK were not well anchored and responded systematically to economic news.Moreover, the model in Gürkaynak, Sack, and Swanson (2005) shows that allof these empirical results are consistent with financial markets expecting somedegree of pass-through (or some chance of pass-through) from short-term inflationto the long-term inflation outlook. In the U.S., the Federal Reserve has no specifictarget for steady-state inflation, only a mandate for “price stability,” which the

28. The Economist’s analysis of the Bank of England’s move, rather than being idiosyncratic, wasechoed throughout the British press at the time. For example, The Financial Times reported the dayafter the move that: “Mr. Kenneth Clarke, the chancellor, boosted his credibility,” that “the Bank ofEngland’s reputation was also enhanced,” and that “the clear message is that the Bank of England hasmuch more independence in setting monetary policy than at any time in its history” (The FinancialTimes, 1994).


Fed has interpreted broadly. In the pre-1997 UK, although there was an officialnumerical inflation target, that target appears to have not been credible due to lackof central bank independence. By contrast, in the UK since independence in 1998,far-ahead forward inflation compensation has been quite stable and insensitive toeconomic news, consistent with the view that a credible inflation target helps toanchor private sector views about the distribution of long-run inflation outcomes.

If this hypothesis is correct, then we might expect to see the effects of acredible inflation target manifesting itself in measures of inflation expectationsother than bond yields and in inflation targeting countries other than the UK. Inthis section, we consider such additional supporting evidence. In particular, weshow that lower-frequency, survey-based measures of inflation expectations in theU.S. and UK corroborate our findings above, and that far-ahead forward inflationcompensation in Sweden, another inflation targeting country with a relativelylong history of inflation-indexed bonds, also appears well-anchored. Finally, wediscuss the robustness of our results to possible time-variation in term premia.

6.1. Disagreement among Professional Forecasters

Cross-country comparisons using survey data are inevitably fraught with diffi-culties, due to differences in the sampling methodology and other idiosyncrasies.Nevertheless, it is useful to gauge the degree of consensus or disagreement amongprofessional forecasters regarding the longer-run inflation outlook for the UnitedKingdom and the United States.

For more than a decade, the Federal Reserve Bank of Philadelphia hasconducted a quarterly Survey of Professional Forecasters (SPF) that includesa projection of the 10-year-average inflation rate for the U.S., and measures ofdispersion across forecasters can be constructed using the entire set of individualprojections. Since 1997, the Bank of England’s Inflation Report has also includeda summary of its latest survey of external forecasters (SEF). Each Bank of Englandsurvey elicited inflation projections based on the price index that was referencedin the Bank’s inflation target; that is, the RPIX in surveys through the last quarterof 2003, and the CPI in surveys since 2004. Moreover, these projections were pro-vided for the annualized one-quarter inflation rate at forecast horizons up to ninequarters ahead—the horizon at which the Bank generally intends that inflationwill be stabilized at the target value. Although individual survey responses arenot reported, the cross-sectional dispersion in forecasters’ inflation projections isevident from the histogram published in each Inflation Report.

Absent any other considerations, one might anticipate that a comparisonof these two surveys would reveal greater dispersion in nine-quarter-ahead UKinflation projections than for 10-year-average U.S. inflation projections. Indeed,even with a transparent and credible inflation target, some shocks to the UKeconomy might well have inflationary consequences lasting longer than two years;


hence, substantial uncertainty—and a corresponding degree of dispersion acrossforecasters—would be evident at a nine-quarter forecast horizon. And if U.S.long-run inflation expectations were firmly anchored, one might anticipate that thecurrent state of the economy would have only modest implications for the averageinflation rate over the subsequent decade, implying that professional forecasters’projections at that horizon would be concentrated in a fairly narrow range.

Despite these limitations, the survey evidence confirms that inflation expec-tations of professional forecasters have been anchored quite firmly in the UnitedKingdom. As depicted in the left panel of Figure 6, about two-thirds of the par-ticipants in the SEF in 2001:Q4 were projecting that UK inflation nine quarterslater would fall within 0.1% of the Bank of England’s inflation target of 2.5%,and nearly all of the participants expected that inflation would fall within 0.4%of the Bank’s target.

The bottom panel of this figure suggests that the Bank of England was rea-sonably successful in communicating about the switch in its inflation measure inlate 2003; that is, by 2004:Q1, the SEF respondents’ nine-quarter-ahead inflationprojections for the HICP were concentrated at the new inflation target of 2%, withrelatively little change in the cross-sectional distribution.

The survey evidence also highlights the extent to which long-run inflationexpectations have not been completely anchored in the United States. Only one-third of the respondents in the 2001:Q4 SPF were projecting the 10-year-averageU.S. inflation rate at roughly the modal value of 2.5%, and the projections ofother respondents were distributed almost uniformly over an interval from 1.5%to 3.2%. Moreover, as of 2004:Q1, the distribution of SPF inflation projectionswas even more dispersed than in late 2001, showing greater heterogeneity ofexpectations of inflation at long horizons.29

6.2. Swedish Bond Yield Data

In Table 5, we report results for regression equation (2) applied to Sweden.Like the UK, Sweden has been an official inflation targeter throughout muchof the 1990s: After abandoning its currency peg in late 1992, the Riksbank (theSwedish central bank) announced in January 1993 that it would adopt an inflationtargeting framework with an official target of 2% that would become effectivebeginning in January 1995 (thus, there is some ambiguity about exactly whatdate should be regarded as the beginning of the inflation targeting regime). Our

29. Note that dispersion of mean beliefs (heterogeneity) and mean dispersion of beliefs (uncer-tainty) are two separate objects. Credible inflation targets are supposed to affect both distributions.The distribution of possible inflation expectations in the far future should be stable and agents shouldhave similar beliefs, centered on the inflation target. The evidence presented here shows that the dis-persion of beliefs across agents corroborates our findings based on the average perceived distributionof possible inflation outcomes reflected in far-ahead forward inflation compensation.


Figure 6. Disagreement in medium-to-long-run inflation projections for the United Kingdom andthe United States. This figure depicts the cross-sectional dispersion in professional forecasters’medium-to-long-run inflation projections in 2001:Q4 (top) and in 2004:Q1 (bottom). The solid barsrepresent the histogram of nine-quarter-ahead projections for the UK RPIX in 2001:Q4 and for theUK of HICP in 2004:Q1, taken from the Bank of England’s quarterly survey of external forecasters.The hatched bars represent the histogram of projections in each period for the 10-year-average U.S.CPI inflation rate, taken from the Federal Reserve Bank of Philadelphia’s Survey of ProfessionalForecasters. In each case, all of the professional forecasters’ projections fell within the range of1.5–3.2%.


Table 5. Swedish forward rate responses to economic news (1996–2005).


Consumer Price Index 1.94∗ 0.85(2.55) (1.13)

Core Consumer Price Index 2.72∗∗ −0.33(4.26) (−0.37)

Real GDP (preliminary) 0.79 0.29(1.17) (0.45)

Industrial Production −0.14 −0.67(−0.24) (−1.55)

Producer Price Index 0.63 −0.24(0.83) (−0.76)

−0.49 0.65Retail Sales (−0.72) (1.21)

−0.26 −0.04Unemployment (−0.67) (−0.11)

Monetary Policy 0.72∗∗ 0.23(3.62) (1.48)

# Observations 514 514R2 .07 .01Joint test p-value .0000∗∗ .420

Notes: Sample: May 1996–Dec 2005 at daily frequency on the dates of macroeconomic and monetary policyannouncements. Heteroskedasticity-consistent t-statistics reported in parentheses. ∗Significant at 5%; ∗∗Significant at1%. Regressions also include a constant, a Y2K dummy that takes on the value 1 on the first business day of 2000, and ayear-end dummy that takes on the value 1 on the first business day of any year (coefficients not reported). Macroeconomicdata release surprises are normalized by their standard deviations, so that coefficients represent a basis point per standarddeviation response. Monetary policy surprises are in basis points, so that those coefficients represent a basis point perbasis point response. Inflation compensation is the difference between nominal and real rates. Joint test p-value is forhypothesis that all eight coefficients (other than the constant and dummy variables) are zero. See text for details.

inflation-indexed bond yield data for Sweden begin in May 1996, so we beginour analysis of Sweden with that date, which has the added advantage of givingthe Riksbank a few years to gain operational experience within the new floatingexchange rate regime and to establish some degree of credibility with respect tothe inflation target (see, e.g., Berg and Grottheim 1997).

As can be seen in Table 5, the results for Sweden are very similar to thosefor the United Kingdom after the Bank of England gained independence, and arestrikingly different from those for the United States or for the pre-independencesample period for the United Kingdom. Although the regression R2 for the shortrate in Sweden is lower than for the U.S. or UK (perhaps because our data on majorSwedish economic announcements is less comprehensive than our data for theU.S. and UK), Swedish short-term interest rates nonetheless respond with a highdegree of statistical significance to three major macroeconomic and monetarypolicy announcements, and the data overwhelmingly reject the hypothesis thatthe one-year rate in Sweden is unrelated to these announcements. Yet far-aheadforward nominal interest rates and inflation compensation in Sweden show no


Figure 7. Daily changes in Swedish far-ahead forward inflation compensation. The left paneldenotes quarterly observations of the percentage point surprise in the advance release of SwedishGDP (normalized by the standard deviation of these surprises) and the corresponding 1-day changein the Swedish 9-year-ahead 1-year forward rate of inflation compensation (in basis points); the rightpanel provides similar information for surprises in the monthly release of the Swedish core consumerprice index. In each panel, the solid line indicates the least-squares regression line.

statistically significant relationship to any of these releases, and the jointhypothesis that the coefficients on all of these releases are zero is not rejected.

The Swedish results are shown graphically in Figure 7. Just as in the morerecent data sample for the United Kingdom, there is no systematic relationshipbetween economic news and far-ahead forward inflation compensation, providingfurther support for the view that an explicit long-run inflation target is helpful inanchoring long-run inflation expectations.

6.3. The Role of Risk Premia

In interpreting and discussing our results to this point, we have emphasized finan-cial market perceptions of the distribution of long-term inflation outcomes. Analternative explanation for our bond yield regression results is one that relies noton changes in the physical or objective distribution of long-term inflation out-comes, but rather on the prices that financial markets place on those risks. Inthis section, we turn to the possible role of time-varying risk premia in explain-ing our bond yield regression findings. Note, however, that our survey-basedmeasures of longer-term inflation expectations in Section 5.1 are in principle com-pletely free of risk premia, providing an additional check on our results and theirinterpretation.

It should be emphasized at the outset that none of our regression results andtheir interpretation require the expectations hypothesis (EH) of the term structure


to hold. According to the EH, long-term bond yields equal the expected return torolling over a series of short-term bonds over the same horizon, plus a possiblynonzero term premium that is constant over time. Although some authors havefound some support for the EH in the data (e.g., Bekaert, Hodrick, and Marshall2001), a number of prominent studies (Fama and Bliss 1987; Campbell and Shiller,1991) have documented strong violations of the EH over a wide variety of samplesand securities, suggesting that the risk, term, liquidity, and/or other premia (oftencollectively referred to as “risk premia”) embedded in long-term bond yields mayin fact vary substantially over time.

However, even if the EH is violated, our results and their interpretationare essentially unaffected so long as those risk premia vary primarily at lowerfrequencies—monthly, annual, or business-cycle frequencies, say. In that case,the change in bond yields on any given day effectively differences out the riskpremium on that day and leaves changes in market expectations as the primarydriver of the 1-day change in yields. Although there is no reason a priori to thinkthat risk premia should vary only at lower frequencies, the predictors of excessreturns on bonds found by these studies in fact do have this feature, in particu-lar, that risk premia vary primarily at business-cycle frequencies (Cochrane andPiazzesi 2005).

Most importantly, however, in order for changes in risk premia to explain ourfindings, one would have to explain why these premia would vary systematicallyin precisely the way that we estimate. For example, it would be a challengeto explain why risk premia might increase in response to positive news aboutoutput and employment and decrease in response to negative news about thesevariables, when a consumption-CAPM model and the results of Cochrane andPiazzesi (2005) and Piazzesi and Swanson (2008) would predict just the opposite.Moreover, it seems difficult to explain why risk premia might respond to economicnews in the U.S. and in the pre-1997 UK, but not in Sweden or in the post-independence UK. The Swedish bond market in particular is much smaller thanthe U.S. market, so one might think that liquidity or other risk premia would, ifanything, be even more of an issue in that country than in the U.S.

None of this is meant to imply that risk premia have been unimportant overour sample—indeed, one possible explanation for the volatility in the forwardinflation compensation series in Figures 1 and 2 is precisely changes in riskpremia over time. It is just that volatility in risk premia by itself is not sufficientto explain all of our results—one would have to explain why those risk premiamove so systematically in response to news in just the way that a pass-throughfrom short-term inflation to longer-term inflation would suggest.

We are sympathetic to the view that changes in the mean expected long-runinflation rate may not be responsible for our findings so much as changes in theskewness or other moments of that distribution, but this idea is in fact entirely con-sistent with our preferred interpretation—namely, that inflation targeting helps to


anchor market perceptions of the entire distribution of future long-run inflationoutcomes.

7. Conclusions

Does inflation targeting help to anchor private sector perceptions of the futuredistribution of long-run inflation outcomes? We find much evidence that it does.In contrast to previous studies using quarterly or even semiannual data, we havepresented evidence from thousands of daily observations of long-term bond yieldresponses to economic news in the U.S., UK, and Sweden that support this conclu-sion. Far-ahead forward inflation compensation in the U.S. and the pre-1997 UKis much more volatile and responds much more significantly to economic newsthan far-ahead forward inflation compensation in the post-1998 UK or Sweden.Longer-term inflation expectations of professional forecasters, taken from surveysin the U.S. and UK, further corroborate these findings.

Our results have potentially important implications for the U.S. Despite thegenerally superb performance of the U.S. economy and U.S. monetary policy inthe 1990s and 2000s, we find that, with respect to long-term interest rates andinflation expectations, the gains realized in the UK and Sweden over this periodhave been even greater. In particular, the Federal Reserve’s informal approach toa long-run inflation objective does not seem to have anchored the private sector’slong-term inflation views to the same extent that we see in the UK and Sweden,both formal inflation targeters. This performance has been all the more remarkablein light of the greatly inferior inflation expectations in the UK and Sweden relativeto the U.S. that existed in the early 1990s.

Although the sensitivity we estimate is only a few basis points per announce-ment, it may be economically important for two reasons. First, although the effectof any single monthly announcement is only a few basis points, the effects ofthese announcements must be cumulated across releases and over time—in otherwords, even though any single monthly data release is just one noisy indica-tor of the state of U.S. economy, the information content of those releases islarger when we consider them jointly and over time. Thus, the few basis pointsper announcement that we estimate often adds up, over the course of just a fewmonths, to large and significant changes in long-term interest rates.30 Second, thesensitivity of far-ahead forward inflation compensation to economic news that weestimate is almost as large as the sensitivity of short-term interest rates to these

30. More technically, the variance of a unit root process increases linearly over time. MonthlyU.S. economic activity and inflation are close enough to unit root processes that the variance of thecumulative sum of monthly data releases increases essentially linearly in the near term. Thus theimpact of these announcements on long-term interest rates will have a tendency to cumulate stronglyin the near term with a variance that grows linearly with time in the near term.


announcements. Because we know that short-term interest rates should and dorespond substantially to current output and inflation, the sensitivity of far-aheadforward inflation compensation that we estimate should also be regarded as beingeconomically as well as statistically significant.

There are many interesting ways to follow up on our findings, but two standout. The first is to study the episode of financial market stress and high commodityprice—when it is over—using the framework of this paper. Central bank responseshave been different across countries and it would be very interesting to find outwhether these differences have been reflected in the behavior of forward inflationcompensation. The second very interesting question to study is expected welfaregains for the U.S. from stabilizing long-term inflation expectations via an explicitinflation target.

Although we have not shown in this paper that there any advantages, eitherqualitatively or quantitatively, to the stabilization of long-term inflation expec-tations, existing macroeconomic and finance theory suggests that there shouldbe several—for example, less persistent deviations of inflation from target inthe near term due to firmer anchoring of expectations at the long end, and agreater ability of the central bank to control inflation in the short and mediumrun (Svensson and Woodford 2003; Woodford 2003; Orphanides and Williams2005); less volatile long-term nominal interest rates and lower risk premia onnominal rates that would improve the efficiency of investment decisions (Inger-soll and Ross 1992); and a reduced chance of a 1970s-style “expectations trap”for inflation (Albanesi, Chari, and Christiano 2003) or an imperfect information-driven “inflation scare” (Orphanides and Williams 2005). To the extent that thesebenefits are important in practice as well as in principle, there are reasons tothink that, with a more explicit inflation objective, U.S. monetary policy and eco-nomic performance could be improved even beyond the successes of the past 20years.

Data Appendix

Data on U.S. macroeconomic statistical releases and forecasts were collectedby Money Market Services up through July 2003, when that company mergedwith a larger financial institution. Subsequent to July 2003, the same surveywas produced again by Action Economics. These data can be purchased fromHaver Analytics as part of the “MMS” series of data at 〈http://www.haver.com〉.

For the UK, we also obtained MMS data for the UK from Haver Analytics.For the UK, many of the MMS series are reported both as month-on-month andyear-on-year changes, so this presents an issue in terms of which version of eachstatistic to use in our analysis. If one version contained many more observationsthan the other, then we used the version that had more observations—in every


case, this turned out to be the year-on-year version (for some series, the month-on-month version was not collected or reported by MMS at all). When the numberof observations was similar across the two versions, we used the year-on-yearchanges by default, because those were the versions that were more often availablein general and those are the versions that are reported in the headlines of the Britishpress. In any case, we have verified that whether we use the month-on-month oryear-on-year changes for the statistical releases has essentially no impact on ourresults.

For Sweden, we obtained data on statistical releases and private sector fore-casts from Bloomberg Financial Services. Bloomberg offers forecasts of a greaternumber of Swedish and Euro area statistics than Money Market Services thatwere more readily available to us. Bloomberg also offers these data for theU.S. and UK as well, although they do not go as far back as the MMS data(about 1996 for Bloomberg vs. 1985 for MMS-U.S. and 1993 for MMS-UK),but the two data sources agree very closely when they overlap. As in the UK,when there is a choice between the year-on-year and month-on-month versionsof a statistical release, we favor the year-on-year version unless there are moredata available for the month-on-month version. Also as in the UK, whether weuse the month-on-month or year-on-year versions has very little effect on ourresults.

The exact statistics we use for all of these countries, including MMS andBloomberg mnemonics, are reported in the following tables.

United States

Name MMS/Haver Mnemonic

core Consumer Price Index {L,M}111CPCMreal GDP (advance) {L,M}111GPAAInitial Jobless Claims {L,M}111ICNAPM/ISM Manufacturing {L,M}111PMIFNew Home Sales {L,M}111HNSTNonfarm Payrolls {L,M}111ED

United Kingdom

Name MMS/Haver Mnemonic

Average Earnings {L,M}112AEPYreal GDP (preliminary) {L,M}112GPAYManufacturing Industrial Production {L,M}112MFPYProducer Price Index {L,M}112PPIYRetail Price Index excl Mortgage Interest {L,M}112RPXYRetail Sales {L,M}112RSRY


Sweden

Name Bloomberg Mnemonic

Consumer Price Index swcpyoycore Consumer Price Index swcpundyreal GDP (preliminary) swgdpwyyIndustrial Production swipnsyyProducer Price Index swppiyoyRetail Sales swrsiyoyUnemployment swue

Monetary Policy Surprises

For the U.S., we measure monetary policy surprises using federal funds futures,which provide high-quality, virtually continuous measures of market expectationsfor the federal funds rate. Gürkaynak, Sack, and Swanson (2007) show that,among the many possible financial market instruments that potentially reflectexpectations of monetary policy, federal funds futures are the best predictor offuture policy actions. The federal funds futures contract for a given month settlesat the end of the month based on the average federal funds rate that was realizedover the course of that month. Thus, daily changes in the current-month futuresrate reflect revisions to the market’s expectations for the federal funds rate over theremainder of the month. (Our results are essentially unchanged if we instead usethe change in a 3-month interest rate, such as the 3-month Treasury bill rate, onthe days of monetary policy announcements.) As explained in Kuttner (2001) andGürkaynak, Sack, & Swanson (2007), the change in the current month’s contractrate on the day of a Federal Open Market Committee (FOMC) announcementcan be scaled up to account for the timing of the announcement within the monthto provide a measure of the surprise component of the FOMC decision.31 Wecompute the surprise component associated with every FOMC meeting and inter-meeting policy action by the FOMC over our sample.32

For the United Kingdom, we do not have futures data for the policy rate of theBank of England, so we measure monetary policy surprises using the change inthe 3-month government bill rate (obtained from the Bank of England’s Web site)on the days of Bank of England monetary policy announcements. The change in

31. In order to avoid very large scale factors, if the monetary policy announcement occurs in the lastseven days of the month, we use the next-month contract rate instead of scaling up the current-monthcontract rate.32. There is one exception in that we exclude the intermeeting 50-basis point easing on 17 Septem-ber 2001 because financial markets were closed for several days prior to that action and because thateasing was a response to a large exogenous shock to the U.S. economy, and we would have difficultydisentangling the effect of the monetary policy action from the effect of the shock itself on financialmarkets that day.


the 3-month rate on these days reflects changes in financial market expectationsabout the current and future course of monetary policy over the subsequent 3months. Although this is not the same as the shorter horizon one would obtainfrom a very near-term futures contract, it is nonetheless an excellent measure ofthe change in the near-term monetary policy environment. When the 1-day changein the government bill rate isn’t available (this is the case for many observationsin the pre-1997 period), we use the 2-day change in the spot 3-month sterlingLondon Interbank Offer Rate (LIBOR). (The LIBOR rate is quoted at around11 a.m. London time and we do not always know at what time the monetarypolicy announcement was made to the public, so we must use a two-day windowto be sure of bracketing the announcement.) Prior to May 1997, the only monetarypolicy announcements that we have are actual changes in the policy rate by theBank of England, the dates of which we obtained from the Bank of England’sWeb site. After May 1997, there are well-defined announcement dates in the eventof no change in policy rate as well. For comparability to the pre-1997 period, wecontinue to focus only on changes in the policy rate from 1998–2005, but ourresults are very similar if we use all monetary policy announcements (includingthose with no rate change) in the post-1998 period.

For Sweden, we also do not have futures data on the monetary policy instru-ment and use the change in the 3-month Swedish Government Bill rate on thedays of Riksbank monetary policy announcements, which we obtained from theSwedish Riksbank’s Web site.

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